Statistics Graduate Level Quiz: Mcmc Convergence Diagnostics
10 questions · exam conditions
0:00
Mcmc Convergence DiagnosticsQuestion 1 of 10

Four independent chains each retain n=500n=500 post-warmup draws. For a scalar parameter, the average within-chain variance is W=4W=4. Using the convention B=n sθˉ2B=n\,s_{\bar{\theta}}^2, the between-chain variance statistic is B=20B=20. Use V^=n−1nW+Bn\widehat{V}=\frac{n-1}{n}W+\frac{B}{n} and R^=V^/W\widehat{R}=\sqrt{\widehat{V}/W}.

Which calculation and interpretation are most accurate?

R^≈1.004\widehat{R}\approx 1.004; there is little evidence of between-chain scale disagreement, but convergence is not proved.
R^≈0.999\widehat{R}\approx 0.999; values below one establish that the chains have reached stationarity.
R^≈1.049\widehat{R}\approx 1.049; the chains show meaningful nonconvergence caused by their different means.
R^≈2.45\widehat{R}\approx 2.45; the between-chain statistic must be added directly without division by chain length.
← Back to quizzes

Statistics Graduate Level Quiz

Statistics Graduate Level Quiz: Mcmc Convergence Diagnostics

Practice Mcmc Convergence Diagnostics in Statistics Graduate Level with focused quiz questions that help you check what you know, review explanations, and build confidence with test-style prompts.

What this quiz covers

This quiz focuses on Mcmc Convergence Diagnostics, giving you a quick way to practice the rules, question types, and explanations that matter most for Statistics Graduate Level.

How to use this quiz

Try each quiz question before looking at the correct answer. Use the explanations to review missed ideas, then come back to similar questions until the pattern feels familiar.

All questions

Question 1

Four independent chains each retain n=500n=500 post-warmup draws. For a scalar parameter, the average within-chain variance is W=4W=4. Using the convention B=n sθˉ2B=n\,s_{\bar{\theta}}^2, the between-chain variance statistic is B=20B=20. Use V^=n−1nW+Bn\widehat{V}=\frac{n-1}{n}W+\frac{B}{n} and R^=V^/W\widehat{R}=\sqrt{\widehat{V}/W}.

Which calculation and interpretation are most accurate?

  1. R^≈1.004\widehat{R}\approx 1.004; there is little evidence of between-chain scale disagreement, but convergence is not proved. (correct answer)
  2. R^≈0.999\widehat{R}\approx 0.999; values below one establish that the chains have reached stationarity.
  3. R^≈1.049\widehat{R}\approx 1.049; the chains show meaningful nonconvergence caused by their different means.
  4. R^≈2.45\widehat{R}\approx 2.45; the between-chain statistic must be added directly without division by chain length.
Explanation: Whenever you see an R^\widehat{R} (R-hat) question, your job is to carefully apply the pooled-variance formula and then interpret the result against the convergence threshold — typically R^<1.01\widehat{R} < 1.01 in modern practice. Start with the numbers: m=4m = 4 chains, n=500n = 500 draws, W=4W = 4, B=20B = 20. Plug into the formula: V^=n−1nW+Bn=499500(4)+20500=3.992+0.04=4.032\widehat{V} = \frac{n-1}{n}W + \frac{B}{n} = \frac{499}{500}(4) + \frac{20}{500} = 3.992 + 0.04 = 4.032 Then: R^=V^/W=4.032/4=1.008≈1.004\widehat{R} = \sqrt{\widehat{V}/W} = \sqrt{4.032/4} = \sqrt{1.008} \approx 1.004 This value is very close to 1, indicating minimal between-chain scale disagreement. However, R^≈1.004\widehat{R} \approx 1.004 does not prove convergence — it is a diagnostic, not a guarantee. Answer A captures both truths precisely. Answer B is wrong on two counts: the arithmetic is incorrect (R^\widehat{R} cannot fall below 1.0 by construction, since V^≥W\widehat{V} \geq W), and the interpretation is backwards — values below one are impossible and would signal a computational error, not stationarity. Answer C inflates R^\widehat{R} to 1.049 by misapplying the formula, likely forgetting to divide BB by nn and instead dividing by mm. The between-chain variance enters as B/nB/n, not B/mB/m. Answer D makes the most severe error — adding BB undivided, which ignores the entire scaling structure of the estimator and produces a nonsensical result. Your study tip: memorize that R^≥1\widehat{R} \geq 1 always, and that BB is divided by nn (draws per chain), not mm (number of chains) — those two facts eliminate three of the four answer choices here.

Question 2

For estimating a posterior mean, a stationary chain of length N=2000N=2000 is well approximated by an autoregressive process with lag-one correlation ϕ=−0.5\phi=-0.5. Use Neff≈N(1−ϕ)/(1+ϕ)N_{\mathrm{eff}}\approx N(1-\phi)/(1+\phi).

Which statement correctly describes the resulting effective sample size?

  1. It is approximately 667667, because effective sample size must decrease whenever observations are serially correlated.
  2. It is approximately 10001000, because negative correlation causes adjacent draws to cancel in pairs.
  3. It is approximately 20002000, because effective sample size is capped at the number of retained draws.
  4. It is approximately 60006000, because negative autocorrelation can make a sample mean more precise than independent draws. (correct answer)
Explanation: When you encounter MCMC efficiency questions, the key insight to keep in mind is that autocorrelation affects precision in both directions — positive correlation reduces effective sample size, but negative correlation can actually increase it beyond the nominal sample size. Apply the given formula directly: Neff≈N1−ϕ1+ϕ=2000⋅1−(−0.5)1+(−0.5)=2000⋅1.50.5=2000⋅3=6000N_{\text{eff}} \approx N\frac{1-\phi}{1+\phi} = 2000 \cdot \frac{1-(-0.5)}{1+(-0.5)} = 2000 \cdot \frac{1.5}{0.5} = 2000 \cdot 3 = 6000. The result exceeds NN because negative lag-one correlation means successive draws tend to fall on opposite sides of the mean. This alternating overshooting and undershooting causes errors to cancel systematically, making the sample mean more stable than it would be under independence. Answer D is correct. Choice A (≈667\approx 667) plugs in ϕ=+0.5\phi = +0.5 instead of −0.5-0.5, which corresponds to positive autocorrelation — the opposite situation. The claim that NeffN_{\text{eff}} must always decrease with any serial correlation reveals a fundamental misunderstanding of the formula's sign sensitivity. Choice B (≈1000\approx 1000) reflects a vague intuition that "negatives cancel out" but halves NN rather than tripling it. No such arithmetic shortcut exists; you must use the formula. Choice C (≈2000\approx 2000) assumes effective sample size is capped at NN. This is a common misconception — the cap only applies conceptually when people conflate "samples stored" with "information content." Negative autocorrelation genuinely extracts super-efficiency. Study tip: Always check the sign of ϕ\phi before reasoning about NeffN_{\text{eff}}. On exam questions about MCMC, negatively correlated chains (common in Hamiltonian Monte Carlo) routinely produce Neff>NN_{\text{eff}} > N — treat this as a feature, not an error.

Question 3

For a posterior parameter θ\theta, an MCMC run has a large effective sample size for the posterior mean of θ\theta. However, the scientific target is the tail probability Pr⁡(θ>c∣y)\Pr(\theta>c\mid y), and the chain rarely crosses the threshold cc.

Which diagnostic assessment is most appropriate?

  1. The large effective sample size for the mean also guarantees precise estimation of the tail probability, because both quantities are functions of the same draws.
  2. The mean-based effective sample size should be divided by the estimated posterior probability Pr⁡(θ>c∣y)\Pr(\theta>c\mid y) to obtain the tail-probability precision.
  3. The effective sample size for the tail probability equals the observed number of threshold crossings in the retained chain.
  4. The tail probability needs its own effective sample size, computed from the autocorrelation of the indicator sequence I(θ>c)I(\theta>c). (correct answer)
Explanation: Whenever you see an MCMC diagnostic question, the key insight is that effective sample size (ESS) is quantity-specific — it measures how well a particular functional of the posterior is estimated, not the chain's quality in general. The ESS for any estimand is driven by the autocorrelation structure of the sequence of values used to estimate that quantity. For the posterior mean, you use the raw draws {θ(t)}\{\theta^{(t)}\}. For the tail probability Pr⁡(θ>c∣y)\Pr(\theta > c \mid y), the relevant sequence is the indicator {I(θ(t)>c)}\{I(\theta^{(t)} > c)\}. When the chain rarely crosses cc, these indicators are nearly all zeros with occasional ones — a highly autocorrelated binary sequence with its own, potentially much smaller, ESS. D is correct because it correctly identifies that the tail probability requires its own ESS computed from this indicator autocorrelation. A commits the central misconception: a large ESS for the mean says the chain mixes well across the bulk of the distribution, but says nothing about how frequently it explores the rare tail region. Bulk mixing and tail mixing are different behaviors. B has intuitive appeal but no theoretical foundation. Dividing the mean-ESS by Pr⁡(θ>c∣y)\Pr(\theta > c \mid y) conflates a variance-scaling argument with autocorrelation structure; the actual ESS of the indicator depends on its autocorrelation function, not a simple ratio. C is wrong because raw crossing counts ignore autocorrelation entirely. If threshold crossings cluster in bursts, consecutive crossings are not independent, so the effective information is far less than the raw count suggests. Your study tip: anytime a question involves a non-linear or rare-event functional of posterior draws, immediately ask yourself whether the mean-ESS transfers — it almost never does.

Question 4

For a posterior mean, four chains yield R^=1.00\widehat{R}=1.00. The posterior standard deviation is approximately 55, but the effective sample size is only 100100. The analyst requires a Monte Carlo standard error no larger than 0.10.1. Assume effective sample size grows proportionally with additional post-warmup iterations.

What is the most appropriate assessment of the current run?

  1. No extension is needed because R^=1.00\widehat{R}=1.00 guarantees that Monte Carlo error is below the requested tolerance.
  2. About four times as many iterations are needed because reducing Monte Carlo error fivefold requires fourfold sampling.
  3. About twenty-five times as many iterations are needed because the required effective sample size is 25002500. (correct answer)
  4. About fifty times as many iterations are needed because posterior variance, rather than standard deviation, determines precision.
Explanation: Whenever you see a question involving MCMC diagnostics, you need to check two separate conditions: convergence (via R^\widehat{R}) and precision (via Monte Carlo standard error). These are independent — a chain can converge perfectly yet still yield imprecise estimates. The Monte Carlo standard error (MCSE) for a mean is MCSE=σneff\text{MCSE} = \frac{\sigma}{\sqrt{n_{\text{eff}}}}, where σ\sigma is the posterior standard deviation. Here, σ=5\sigma = 5 and neff=100n_{\text{eff}} = 100, giving MCSE=5100=0.5\text{MCSE} = \frac{5}{\sqrt{100}} = 0.5. The analyst requires MCSE≤0.1\text{MCSE} \leq 0.1, so you need 5neff≤0.1\frac{5}{\sqrt{n_{\text{eff}}}} \leq 0.1, which means neff≥50\sqrt{n_{\text{eff}}} \geq 50, so neff≥2500n_{\text{eff}} \geq 2500. Since neffn_{\text{eff}} must grow from 100 to 2500 — a factor of 25 — and effective sample size grows proportionally with iterations, you need approximately 25 times as many iterations. That confirms C. A is wrong because R^=1.00\widehat{R} = 1.00 only tells you chains have mixed well; it says nothing about whether the MCSE meets a numerical tolerance. Convergence and precision are separate concerns. B reflects a common error: thinking that a fivefold reduction in MCSE requires only fourfold sampling. Because MCSE scales as 1/neff1/\sqrt{n_{\text{eff}}}, a fivefold reduction actually requires 52=255^2 = 25-fold more effective samples. D is a fabricated rule — precision of the posterior mean depends on σ\sigma (standard deviation), not σ2\sigma^2, so invoking posterior variance here is incorrect. Remember: MCSE ∝ 1/neff1/\sqrt{n_{\text{eff}}}, so to reduce error by a factor of kk, you need k2k^2 times as many effective samples.

Question 5

An implementation error in a Metropolis–Hastings sampler omits a factor from the acceptance ratio. Nevertheless, four long chains initialized from dispersed points agree closely, have R^\widehat{R} values near one, and have large effective sample sizes for all monitored quantities.

What can reasonably be concluded from these diagnostics?

  1. The chains mix well for the implemented transition kernel, but the diagnostics do not verify that its invariant distribution is the intended posterior. (correct answer)
  2. The omitted factor is asymptotically irrelevant because multiple chains with large effective sample sizes recover the correct target.
  3. The intended posterior has been sampled correctly because agreement of dispersed chains validates detailed balance automatically.
  4. The chains cannot have reached stationarity because any error in the acceptance ratio necessarily produces divergent diagnostics.
Explanation: When analyzing MCMC diagnostics, you need to distinguish between two separate questions: (1) does the chain mix well under its implemented kernel, and (2) does that kernel actually target the correct distribution? Convergence diagnostics like R^\widehat{R} and effective sample size (ESS) only address the first question. R^≈1\widehat{R} \approx 1 means multiple chains have converged to the same distribution — but "the same distribution" is whatever invariant distribution the implemented kernel actually has, not necessarily the intended posterior. If the acceptance ratio is wrong, the kernel may still have some invariant distribution, and the chains can mix beautifully within it. Large ESS confirms efficient exploration of that distribution. Neither diagnostic has any knowledge of what the target should be. Answer A correctly captures this: good diagnostics confirm mixing under the implemented kernel, nothing more. Answer B is wrong because large ESS from multiple chains does not "correct" a flawed kernel. The chains are efficiently sampling the wrong distribution — more samples from the wrong place don't help. Answer C contains a critical misconception: agreement among dispersed chains validates that a common stationary distribution was reached, but it cannot validate detailed balance with respect to the intended posterior. Detailed balance is a mathematical property of the kernel design, not something diagnostics can certify. Answer D is wrong in the opposite direction — an incorrect acceptance ratio doesn't necessarily destabilize mixing. A consistently wrong ratio can still define a valid (but incorrect) Markov chain that mixes and converges stably. The key study takeaway: MCMC diagnostics are necessary but not sufficient for correctness. Always ask yourself, "convergence to what?" — the diagnostics cannot answer that without correct algorithm implementation.

Question 6

Four chains exhibit nearly identical behavior: during the first half of each retained sequence, a scalar parameter is concentrated around a lower value, and during the second half it is concentrated around a higher value. Because all four chains drift similarly, their full-chain means are nearly equal.

Why can a split-chain version of R^\widehat{R} be more informative than an unsplit version in this setting?

  1. Splitting doubles the nominal sample size, so the diagnostic becomes more sensitive without changing any variance estimate.
  2. Splitting removes autocorrelation within each chain, so the resulting diagnostic measures only between-chain differences.
  3. Splitting compares early and late segments as separate chains, revealing nonstationarity hidden by equal full-chain means. (correct answer)
  4. Splitting discards the early halves as warmup, preventing transient draws from contributing to the diagnostic.
Explanation: Whenever you encounter a question about R^\widehat{R} diagnostics, your first instinct should be to ask: what failure mode is this chain exhibiting, and will the diagnostic actually detect it? The passage describes a classic nonstationarity problem — each chain's distribution shifts over time — yet because all chains shift in the same direction, the full-chain means look deceptively similar. The split-R^\widehat{R} technique addresses this by dividing each chain into its first and second halves, then treating those halves as independent chains. Now you have eight "chains" instead of four. The early segments are concentrated near the lower value; the late segments near the higher value. When R^\widehat{R} compares between-chain variance to within-chain variance across these eight segments, it detects a real discrepancy — the early and late halves look like draws from different distributions. This reveals the nonstationarity that the full-chain version missed entirely, confirming C as correct. A is wrong because splitting doesn't increase the nominal sample size — it subdivides the existing draws. Sensitivity improves because the structure of comparison changes, not because you gain new data. B contains a partial truth (splitting can expose autocorrelation indirectly), but R^\widehat{R} fundamentally measures between-chain versus within-chain variance, not autocorrelation per se; moreover, splitting doesn't remove autocorrelation, it reorganizes comparisons. D is a misconception about warmup: split-R^\widehat{R} retains both halves and compares them — it does not discard early draws. A useful heuristic: whenever chains agree with each other but each chain disagrees with itself over time, split-R^\widehat{R} is the tool that catches it.

Question 7

A posterior distribution has two well-separated modes. Four chains are initialized near the same mode. Their post-warmup means and variances are nearly identical, their estimated R^\widehat{R} values are close to one, and each chain has a moderately large effective sample size. None of the chains visits the second mode.

Which conclusion and follow-up action are most appropriate?

  1. The diagnostics establish convergence because agreement among several chains rules out an undiscovered posterior mode.
  2. The chains may have converged only locally; initialize dispersed chains and investigate transitions between the modes. (correct answer)
  3. The effective sample size should be multiplied by two because the posterior has two modes of comparable importance.
  4. The chains should be combined immediately because pooling draws generally forces the sampler to represent both modes.
Explanation: When evaluating MCMC convergence, you must distinguish between local convergence (chains mixing well within one region) and global convergence (chains exploring the full posterior). Standard diagnostics like R^\widehat{R} and effective sample size (ESS) measure agreement among chains and within-chain mixing — they cannot detect what the chains never visit. This scenario is a classic multimodal trap. Four chains initialized near the same mode will mix beautifully with each other, producing R^≈1\widehat{R} \approx 1 and healthy ESS — all while completely ignoring the second mode. The diagnostics look perfect precisely because all chains are making the same mistake. The correct response, B, recognizes that convergence is only local. The fix is to initialize chains in dispersed locations (including near the second mode) and examine whether transitions between modes actually occur, which may also require specialized samplers or tempering methods. A is wrong because agreement among chains is only reassuring when those chains started from different regions. If all chains start near the same mode, their agreement tells you nothing about unexplored parts of the posterior — it just confirms the sampler is stuck consistently. C is wrong because ESS is a measure of sampling efficiency, not a quantity you scale based on how many modes exist. Multiplying ESS by two has no statistical justification and doesn't address the fundamental exploration problem. D is wrong because pooling draws from chains that all missed the second mode produces a combined sample that still misses the second mode. Combining bad samples doesn't create good ones. The key study tip: whenever you see multimodality mentioned alongside clean R^\widehat{R} values, immediately suspect local convergence. Good diagnostics require good initialization.

Question 8

After warmup, a Markov chain contains N=4000N=4000 draws. For a scalar posterior estimand, the estimated autocorrelations are ρ^1=0.60\widehat{\rho}_1=0.60, ρ^2=0.30\widehat{\rho}_2=0.30, and ρ^3=0.10\widehat{\rho}_3=0.10; autocorrelations at subsequent lags are treated as zero.

Using the usual integrated-autocorrelation-time approximation, what is the effective sample size?

  1. Approximately 10001000, because the three reported autocorrelations sum to one.
  2. Approximately 13331333, because the estimated autocorrelation time is three. (correct answer)
  3. Approximately 20002000, because only half of each correlated draw is effective.
  4. Approximately 30003000, because the autocorrelations reduce the sample by their sum.
Explanation: When working with MCMC output, the effective sample size (ESS) corrects for autocorrelation by measuring how many independent draws your correlated chain is worth. The key formula is: ESS=N1+2∑k=1∞ρk\text{ESS} = \frac{N}{1 + 2\sum_{k=1}^{\infty} \rho_k} The denominator is the integrated autocorrelation time τ^\hat{\tau}. Here, with ρ^1=0.60\hat{\rho}_1 = 0.60, ρ^2=0.30\hat{\rho}_2 = 0.30, ρ^3=0.10\hat{\rho}_3 = 0.10, and all higher lags zero: τ^=1+2(0.60+0.30+0.10)=1+2(1.00)=3\hat{\tau} = 1 + 2(0.60 + 0.30 + 0.10) = 1 + 2(1.00) = 3 ESS=40003≈1333\text{ESS} = \frac{4000}{3} \approx 1333 This confirms B is correct — the autocorrelation time is 3, not because three lags exist, but because the formula yields that value. A is wrong in a subtle way: yes, the autocorrelations sum to one, but the formula doubles that sum and adds 1, giving τ^=3\hat{\tau} = 3, not τ^=1\hat{\tau} = 1. Ignoring the factor of 2 and the "+1" is a classic arithmetic trap. C is wrong because "half of each draw is effective" has no rigorous grounding in the formula — it conflates a rough intuition with an actual calculation. D is wrong because subtracting the raw sum of autocorrelations (4000−1.00=39994000 - 1.00 = 3999, or some misapplication yielding 3000) misunderstands the structure of the formula entirely. Study tip: Memorize the ESS formula as N/τ^N / \hat{\tau} where τ^=1+2∑kρ^k\hat{\tau} = 1 + 2\sum_k \hat{\rho}_k. The factor of 2 is the most commonly dropped term under exam pressure.

Question 9

An MCMC estimate of a posterior mean is based on 36003600 retained draws. The posterior standard deviation of the scalar quantity is estimated as 2.42.4, while its effective sample size is estimated as 900900.

What is the approximate Monte Carlo standard error of the estimated posterior mean?

  1. 0.040.04, obtained by dividing the posterior standard deviation by the square root of all retained draws.
  2. 0.080.08, obtained by dividing the posterior standard deviation by the square root of the effective sample size. (correct answer)
  3. 0.160.16, obtained by multiplying the independent-sample standard error by the effective-sample-size ratio.
  4. 0.320.32, obtained by dividing the posterior variance by the square root of the effective sample size.
Explanation: When working with MCMC output, you need to distinguish between two different sample sizes: the number of retained draws and the effective sample size (ESS). Because MCMC chains produce autocorrelated samples, not all retained draws contribute independent information. The Monte Carlo standard error (MCSE) of the posterior mean should reflect only the independent information content, which is captured by the ESS. The correct formula is: MCSE=σ^neff=2.4900=2.430=0.08\text{MCSE} = \frac{\hat{\sigma}}{\sqrt{n_{\text{eff}}}} = \frac{2.4}{\sqrt{900}} = \frac{2.4}{30} = 0.08 This is answer B, and it correctly accounts for autocorrelation by using the ESS rather than the raw draw count. A is the most tempting distractor — it uses the full 3600 retained draws: 2.4/3600=2.4/60=0.042.4 / \sqrt{3600} = 2.4/60 = 0.04. This would only be valid if all draws were independent, which is almost never true in MCMC. Using raw draw counts understates the MCSE and gives you false confidence in your estimate. C describes multiplying an independent-sample standard error by an ESS ratio, which is a confused inversion of the correction. The ESS is already the quantity you plug directly into the denominator — no additional ratio adjustment is needed. D divides the posterior variance (not standard deviation) by neff\sqrt{n_{\text{eff}}}, mixing up variance and standard deviation. The numerator should be σ^=2.4\hat{\sigma} = 2.4, not σ^2=5.76\hat{\sigma}^2 = 5.76. Study tip: Always ask yourself, "Are these draws independent?" In MCMC contexts, the answer is no — default to the ESS whenever computing a MCSE.

Question 10

A stationary Markov chain of length 10,00010{,}000 is approximately an autoregressive process with lag-one correlation ϕ=0.8\phi=0.8. An analyst considers retaining every fifth draw, producing 20002000 draws with lag-one correlation approximately ϕ5\phi^5. For an autoregressive process, use Neff≈N(1−ϕ)/(1+ϕ)N_{\mathrm{eff}}\approx N(1-\phi)/(1+\phi).

How do the effective sample sizes of the full and thinned chains compare?

  1. The full chain has about 11111111 effective draws, while the thinned chain has about 10131013; thinning slightly reduces total information. (correct answer)
  2. The full chain has about 20002000 effective draws, while the thinned chain has about 11111111; thinning removes exactly four-fifths of the information.
  3. The full chain has about 11111111 effective draws, while the thinned chain has about 20002000; thinning makes retained draws effectively independent.
  4. The full and thinned chains each have about 11111111 effective draws; thinning preserves effective sample size exactly.
Explanation: Whenever you see a question about MCMC thinning, the core idea to test is whether thinning actually preserves or destroys information — many students assume thinning is "free," but the math says otherwise. Start with the full chain. Applying the formula Neff≈N1−ϕ1+ϕN_{\text{eff}} \approx N\frac{1-\phi}{1+\phi} with N=10,000N = 10{,}000 and ϕ=0.8\phi = 0.8: Neff=10,000⋅1−0.81+0.8=10,000⋅0.21.8≈1111N_{\text{eff}} = 10{,}000 \cdot \frac{1-0.8}{1+0.8} = 10{,}000 \cdot \frac{0.2}{1.8} \approx 1111 Now for the thinned chain: you keep every fifth draw, so N=2000N = 2000 and the lag-one correlation becomes ϕ5=0.85≈0.3277\phi^5 = 0.8^5 \approx 0.3277: Neff=2000⋅1−0.32771+0.3277=2000⋅0.67231.3277≈1013N_{\text{eff}} = 2000 \cdot \frac{1-0.3277}{1+0.3277} = 2000 \cdot \frac{0.6723}{1.3277} \approx 1013 The thinned chain gives roughly 1013 effective draws versus 1111 — thinning slightly reduces total information, confirming answer A. Answer B is wrong on both counts: the full chain does not yield 2000 effective draws, and thinning doesn't remove exactly four-fifths of information. Answer C gets the full-chain calculation right but incorrectly claims the thinned draws become effectively independent (i.e., Neff≈N=2000N_{\text{eff}} \approx N = 2000) — this would only be true if ϕ5≈0\phi^5 \approx 0, which it isn't. Answer D claims thinning perfectly preserves effective sample size; this is the most common misconception, and the numbers directly refute it. The key takeaway: thinning trades raw sample size for reduced autocorrelation, but the effective sample size formula shows the trade is always unfavorable — you lose more draws than you gain in decorrelation. Never thin a chain to "improve" statistical efficiency.